POLICY RESEARCH WORKING PAPER 2157
How Regional Blocs Affect Price data on exports to Brazil
from countries excluded from
Excluded Countries MERCOSUR show that
preferential trading
agreements hurt nonmember
The Price Effects of MERCOSUR countries by compelling them
to reduce their prices to meet
Won Chang competition from suppliers
L. Alan Winters within the regional trading
bloc.
FILE COPY
The World Bank
Development Research Group
Trade
August 1999
Poi (. RvsFAR(CII WORKING, PAPER 2157
Summary findings
The welfare effects of preferential trading agreements are prices of nonmembers' exports to the bloc. T hese can he
most directly linked to changes in trade prices - that is, explained largely by tariff preferences offerec to a
the terms of trade. country's partners.
Chang and Winters use a simple strategic pricing game Focusing on the Brazilian market (by far th largest in
itn segmented markets to measture the effects of MERCOSUR), they show that noninembers' ?xport
MERCOSLJR on the pricing of "nonmenmber" exports to prices to Brazil respond to both most-favorable-nation
the regional trading bloc. Working with detailed data on and preferential tariffs. Preferential tariffs inc uce
unit values and tariffs, they find that the creation of reductions in nonmember export prices.
MERCOStJR is associated with significant declines in the
This paper - a product of Trade, Development Research Group - is part of a larger effort in the group to un derstand the
effects of regional integrationi. (Copies of the paper are available free from the World Bank, 1818 H Street NW, Washington,
DC 20433. Please contact Lili Tabada, room MC3-333, telephone 202-473-6896, fax 202-522-1159, Intcrniet address
Itabada@ worldbank.org. Policy Research Working Papers are also posted on the Web at http://wwNw.worldbanik.org/html/
dec/Publications/Workpapers/lhome.html. The authors may be contacted at wclhang(iworldbank.org or l.a.winters
(( sussex.ac.uk. AUgust 1999. (57 pages)
| The Policy Research W'orkinig Paper Series disseminilates the findings of uwork in progress to enconrage the exchange of idea.: abot
|development issues. Ant objective of the series is to get the findings out quickly, even ifthe presentations are less than fully polish d. The
papers carry the namies of the authors and shozild be cited accordinigly. The findinigs, interpretations, and conclusions expressed in this
paper are entirely those of the authors. They do not necessarily represent the viewt of the W(orld Bank, its Executive Directors or tbe
COU1ntries they represent.
Produced by the Policy Research Dissemination Center
How Regional Blocs Affect Excluded Countries:
The Price Effects of MERCOSUR*
Won Chang f and L. Alan Winters
Keywords: Regional Integration; Terms of Trade; Imperfect Competition; MERCOSUR
JEL classification: F13; F15; C33
t Won Chang is a research student at Columbia University, E-mail: wchang@worldbank.org.
$ L. Alan Winters is Professor of Economics, School of Social Sciences, University of Sussex, Falmer,
BRIGHTON, BN1 9QN, UK. Tel.: +44 (0) 1273 877273; Fax: +44 (0) 1273 673563/678466; E-mail:
L.A.Winters@Sussex.ac.uk; Centre for Economic Policy Research, 90-98, Goswell Road, London, ECIV
7DB, UK; and Centre for Economic Performance, London School of Economics, Houghton Street, London
WC2A 2AE, UK.
* This work was partly conducted while the authors were Consultant and Research Manager in the
Development Research Group of the World Bank. The views expressed in this paper are those of the authors
and should therefore not be attributed to the World Bank or its member governments. The authors are
grateful to Kyle Bagwell, Jagdish Bhagwati, Stephen Cameron, Richard Clarida, Antoni Estevadeordal,
Junichi Goto, Ann Harrison, Ken Leonard, Will Martin, John McLaren, Andrew Newell, Robert Mundell,
Marcelo Olarreaga, Maurice Schiff, Forhad Shilpi, Isidro Soloaga, Anthony Venables and Stan Wellisz for
excellent comments and participation in the seminars at the Inter-American Development Bank, the World
Bank, the US International Trade Commission, the University of Sussex and Columbia University.
1. INTRODUCTION
1.1 Introduction
Preferential Trading Arrangements (PTAs) have now become an integral and
enduring aspect of the multilateral trading regime. Between 1990 and 1997, 87 PTAs were
notified to the WTO, and nearly all signatories of the WTO are currently members of at
least one PTA. Despite such widespread existence, concerns continue about the welfare
impacts of PTAs, especially on excluded countries. The effects of PTAs on the volume
and quantities of trade are studied quite frequently but, as Winters (1997a, b) argues, these
variables are not a reliable guide to welfare effects for non-member countries. The latter
are more directly related to price effects, and of these there are few studies. Indeed, there
is, to our knowledge, no published ex post study of the price effects of a PTA on its trading
partners.
This paper studies one of the most recently formed and controversial customs
unions, MERCOSUR (between Argentina, Brazil, Paraguay, and Uruguay). It examines
the effect that MERCOSUR has had on the prices of its imports from non-members,
assuming that those countries export to two segmented markets, (1) Brazil and (2) rest of
the world, in an imperfectly competitive setting with differentiated products. We
concentrate on the Brazilian import market since it is a large market for imports, by far, the
largest in MERCOSUR and it provides good data over the time period of interest.! We
' Yeats (1998) first raised the question of whether MERCOSUR may be a concern for non-members, since
the most rapidly growing intra-MERCOSUR exports appear to be in products in which members do not have
1
postulate that changes in Brazilian m.f.n. tariff rates led directly to price changes by non-
member firms exporting to Brazil, and that tariff preferences offered to members, e .g.
Argentina, lead to additional 'strategic' price responses within the Brazilian market. We
seek to identify both such responses in commodity-level import data from Brazil and in
export data from its major overseas suppliers.
MERCOSUR nations have made significant tariff adjustments over our sample
period (1989-1996). In addition to unilateral reforrns over 1989-95, they largely abolished
tariffs on imports from partners over 1991-95, as governed by the Treaty of Asunci6n,
1991. MERCOSUR's common external tariff (CET) is based on the Ouro Preto Protocol,
agreed, after much contention, at the end of 1994 and implemented over the following two
years. The different phasing of these adjustments, plus the exceptions to both the CET and
internal free trade-see Olarreaga and Soloaga (1998)-mean that the margins of preference
on internal trade show considerable variation both through time and across commodities.
This helps us to identify their effects empirically.
In the remainder of the paper, Section 1.2 summarizes the literature on the effects
of PTAs on non-members and on identifying price effects empirically. Section 1.3
discusses some stylized facts and descriptive statistics on the major exporters to the
Brazilian market. The formation of MERCOSUR seems likely to have had an immediate
effect on the pricing of non-member exports to the Brazilian market. The Treaty of
Asunci6n cut members' internal tariffs by more than 50% of the m.f.n. rate at the end of
a comparative advantage. Nagarajan (1998) argues instead that intra-regional trade should be compared w ith
extra-regional imports, not extra-regional exports, and that by focusing on the latter, Yeats may exaggerate
the effects of MERCOSUR. Our work is quite different, referring to the prices not the values of trade flows.
2
1991, with the rest of the cut to zero following over the next four years. Intuitively, the
response to such a large discriminatory tariff cut should be for members to increase their
pre-tariff prices, while non-members reduce theirs.
Section 2 briefly presents a model of this process. From this we derive reduced
form estimation equations and a comparative statics exercise (Appendix I) to interpret their
coefficients. The model has two firms, a 'non-member' and a 'member' firm, exporting a
differentiated product to the Brazilian market. The two firms respond to each other's
prices (as well as to their own tariffs, exchange rates, and wages), playing a Bertrand
pricing game within the Brazilian market. We explore the game by examining relative
member and non-member prices in Brazil, and, for certain exporters, the relative prices of
exports to Brazil and to other markets.
Section 3 presents the empirical implementation of the reduced form equations
solved in section 2. It also provides details of MERCOSUR's tariff policy during the
integration period and of the data and their limitations. Section 4 examines the final results
which suggest strongly that m.fn. tariff changes and preferential tariffs both affect supplier
prices significantly, and that MERCOSUR's preferential tariffs caused significant declines,
ceteris paribus, in the prices of non-members' exports to Brazil.
1.2 Brief survey and motivation for the study
One of the major influences on the welfare of any trading economy is its terms of
trade, and thus questions surrounding trade policy should be concerned with this variable.
3
But given its importance in theory this issue is addressed surprisingly rarely in empirical
studies. A seminal contribution was Kreinin (1961) who considered the effects of US
m.f.n. tariff concessions during the post-war years. Kreinin notes that a reduction in US
tariffs would most immediately affect import prices and that only through this medium
would changes in the volume of imports occur. He also shows that US m.f.n. tariff
concessions did indeed lead to considerable changes in foreign export prices.2
By the same token the empirical analysis of the effects of PTAs should be at least
as concerned with price as with volume effects. An elegant but relatively unremarked
theoretical examination of the terms of trade effect of regional integration is given by
Mundell (1964). He elucidates the terms of trade effects in a 3-country model in which
goods are gross substitutes, and in which price changes occur to restore balance of
payments equilibrium after an initial preferential tariff shock occurs. He shows that for a
single tariff change by one member, the preferred exporting partner's terms of trade
unambiguously improve, while the excluded country's deteriorate. The net effect of the
active country's tariff concessions on its own terms of trade is ambiguous, but when two
countries swap preferential concessions, as in a PTA, they collectively improve their terns
of trade vis-a-vis the rest of the world.
More recent studies focusing on PTAs such as Bagwell and Staiger (1998, 1999)
also show that the multilateral negotiations of the GATT and its principles of reciprocity
and non-discrimination foster efficient outcomes which allow governments to escape from
2Kreinin states that "less than a third...of the tariff concessions granted by the US were passed on to the IJS
consumer in the form of reduced import prices, while more than two-thirds.. .accrued to the foreign suppliers
4
a terms of trade driven Prisoners' Dilemma. The authors argue that PTA formation could
enable member countries to exploit greater market power over their terms of trade and
potentially undermine the efficient outcome of multilateral negotiations.
The last result is potentially very significant, for the terms of trade is by far the
most direct way in which PTAs affect the rest of the world (RoW). Precisely paralleling
Kreinin's complaint, the usual empirical approach to assessing the effects of a PTA is to
ask whether, as a result of integration, the RoW's exports to the integrating bloc increase
(which is held to be good) or decrease (bad). Winters (1997a) shows that this is a very
inadequate indicator: first, RoW welfare will be related to its imports not its exports, and
second, in a competitive economy, marginal changes in quantities hardly matter, whereas
changes in the prices of traded goods matter considerably.3 Given that the theoretical
literature focuses so heavily on terms of trade effects, it is surprising that ex-post studies
which examine these variables are so very sparse.
Turning to quantitative studies of the effects of integration, Winters (1997b)
observes that the RoW's terms of trade do figure in a number of ex ante studies (although
frequently with little emphasis), but that no ex post study addresses the issue. Winters and
Chang (forthcoming) started to do so in the case of Spanish accession to the EC, but were
severely hampered by a number of intractable data difficulties. This paper continues our
efforts in a much more satisfactory empirical environment and generates stronger and more
and improved the terms of trade of the exporting nations."
3 Winters also argues that, contrary to the common belief, Kemp and Wan (1976) said nothing about whether
RoW's welfare increases or decreases in the face of a PTA. They showed how it could be kept constant,
completely obviating the need to discuss its determinants.
5
interesting results. Our focus is primarily on how regional schemes affect excluded
countries: specifically, the effect that MERCOSUR has had on the prices of imports in
Brazil since 1991.
A useful empirical literature, on which we build, relies on the micro-foundations of
imperfectly competitive and segmented markets. The 'pass-through' literature attempts to
explain the lack of import price changes following changes in the exchange rate, and the
consequent implication that foreign suppliers' markups change.4 Feenstra (1989) estimates
a markup model for the US markets for motorcycles and trucks and obtains the usefi.l
result that changes in the exchange rate and in tariffs have equal effects on the net price of
imports--the so-called 'symmetry' hypothesis. Feenstra, however, considered only the
rivalry between domestic and imported varieties and so examined only the pass-through cf
the m.f.n. tariff. For the purpose of examining PTAs, however, we have to model the
pricing game that occurs between rival foreign suppliers within a market under
consideration. In imperfectly competitive settings, a firm's pricing depends not only on the
tariff charged on its own product, but also on that charged on its rivals'. If a member-
country firm receives a preferential tariff concession it becomes more competitive in PTA
markets, and non-member firms are likely (although not bound) to reduce their prices in
compensation. With this in mind we move on to present some stylized results and
descriptive statistics.
6
1.3 Stylized results and descriptive statistics
We present three simple calculations of the mean changes in prices (unit values)
since the formation of MERCOSUR5: for various suppliers, the average price of exports to
Brazil relative to those to non-integrating markets (RoW); the prices of exports to Brazil
and RoW in absolute terms; and, using Brazilian data, the relative prices of imports from
members (Argentina) and non-members. To render commodities comparable, the starting
year price has been normalized to be I for each commodity so that we are essentially
measuring price changes. To be precise we estimate and plot the following statistics:
in Figure 1: - In n(s 2 D, i=(1,...,N) and t=(l,...,T),
N j=, Pl90/P2i90
in Figure 2: IN n 5l$i) , i=(1,...,N) and t=(l,...,T),
in Figure 3: IN ,i=(1,...,N) and t=(1,...,T).
N =1 Pl1i'90 /p,i90g
4 Several recent studies analyze incomplete pass-through in the face of exchange rate fluctuations: for
example, theoretical papers by Baldwin (1988), Dornbusch (1987) and Krugman (1987), and cross-sectional
industry empirics by Knetter (1989), Froot and Klemperer (1989) and Schembri (1989).
5 Because no price data are available we have to use unit value data, but since these are available at the 6-
digit level of the Harmonized System (HS-6) which distinguishes 5113 commodities, we can have reasonable
confidence in their accuracy. The 6-digit Harmonized System became the standard classification for trade
and tariff data across countries starting in 1989. Unfortunately, many countries started reporting well after
that date, and there is no other way to obtain data of this level and precision for earlier years.
7
Where the first subscript, I or 2, represents prices paid in Brazil and RoW respectively, the
second, i=l,...,N, the commodity, and the third, t=l,...,T, time, with the beginning year as
base. The bars above the prices indicate that these are pre-tariff prices, and the superscript
$ denotes prices in dollars. We have averaged prices only over the set of commodities for
which we have observations for all years for both markets or suppliers.
Figure 1 presents mean export prices for four major exporters to Brazil and RoW:
the USA (for which 1356 commodities were exported to both markets in all years), Japan
(580), Korea (99), and Argentina (686). The broken lines give the 95% confidence interval
about the means. To infer from Figure 1 an effect of MERCOSUR on prices, we have
implicitly to employ RoW as the 'anti-monde'. On this basis non-members' relative prices
of exports to Brazil declined by approximately 15% between 1991 and 1996.6 Conversely,
for the integrating partner, Argentina, relative pre-tariff prices to Brazil increased. This
latter result is not significantly different from no change, however, possibly because data
on the critical years 1991 and 1992, during which the major shocks occured, are missing.
It is also interesting to see the pattern of the absolute export prices in Figure 2. For
the USA and Korea absolute export prices declined by about 10% following the shock ol-
MERCOSUR, and then began to rise somewhat afterwards. For Japan, absolute dollar
prices to Brazil rose (presumably reflecting the yen's appreciation) but by less than exporn
prices in general.
6 Similar results for USA exports have been obtained using the data provided in Feenstra (1997).
8
Finally, Figure 3 shows relative member/non-member import prices in the Brazilian
market. Argentina's pre-tariff prices rise relative to USA, Korea, and the world as an
aggregate. Japan is different presumably again explained by the appreciating Yen during
the 1990-1995 period.7
These descriptive statistics match our a priori expectations surprisingly well.
Moreover, they refer to significant volumes of international trade. In 1996, for example,
Brazil imports of goods amounted to $56.5 billion: $12.5 billion from the USA (22.2% of
the total), $7.1 billion from Argentina (12.6%), $5 billion from Germany (8.8%), $3.1
billion from Italy (5.4%), and $2.9 billion from Japan (5.1%). Other large suppliers
examined are Korea and Chile, which account for $1.3 and $1.0 billion, (with 2.2 and 1.8%
import share) respectively. At the commodity level the USA has a share of 10% or more of
Brazilian imports in 60% of the HS-6 headings, Argentina in 17%, Germany in 30%, Italy
in 16%, and Japan in 12%. Korea and Chile each have approximately 5% of HS-6
headings which have 10% or greater import share.
2. THE MODEL
2.1 Export Pricing under Imperfect Competition and Segmented Markets
While the pricing figures above are very informative, they are also very crude, and
so we now include a series of controls to model the effects of MERCOSUR more formally.
7 The Yen appreciated by 54% from 144.8 in 1990 to 94.1 Yen/$ in 1995.
9
We use a parsimonious model of export pricing to illustrate the effects we expect to find.
For each good we distinguish two segmented markets, Brazil and the Rest of the Worlcl
(RoW), and two exporting firms, a non-member firm from outside MERCOSUR and a
member firm from inside (always Argentina in our case).8 The firms supply differentiatecl
products9 and maximize profits in their own currency by manipulating duty-paid prices in
their markets (p). They take their input costs, exchange rates and tariffs as given. Costs
(c(x,w)) are homogeneous of degree one in the price of a composite factor, loosely
referred to here as the wage (w). Thus c (x, w) = wc(x), where x is output and c(x) is unil
costs.
The demand for the non-member's differentiated product in Brazil (market 1) is
given by, xI(p1,p1t,Q1,YI), a function of the its own price, p, its major rival's (Argentina)
product price, p*, the aggregate price index, Q, and nominal national income, Y, in Brazil.
The demand for its product in the RoW (market 2) is a function of its own price, the
aggregate price level and national income in RoW, x2(p2,Q2,Y2). We are assuming here
that Argentina is a sufficiently large supplier to the Brazilian market that the non-member
firm's demand may be related to Argentina's prices, but that it is so insignificant in RoW
markets that no separate Argentina price effect will be identifiable.' The non-member
firm's objective function and first order conditions may thus be written:
' We concentrate on the two largest traders of MERCOSUR, Argentina and Brazil because data on Paraguay
and Uruguay are so sparse.
9 We use Arnington's (1969) distinction between a 'good' and 'product'. 'Goods' are distinguished only by
kind whereas 'products' are distinguished by kind and origin of supply.
10
Max [PI XI(PI 'PX l Ql XY ) + e2 P2X2 (P2,I Q2, Y2 ) - Cl (XlI)W - C2 (X2 )W(l)
P,,P2 T2
with F.O.C.s
plllw+ W Clx(XI(P(. PiIQQIY D)) =0 i=- P' (la)
P21 + C W]- 2e C2x (X2 (P2 Q22)) = ° 772p = &2 P2 (lb)
where, in addition to the variables already defined, x1, and t2 are the ad-valorem tariff
factors (I+t) charged by Brazil and RoW, and e, and e2, the supplier countries' currency
prices of a Brazilian REAL and RoW currency. Note that price elasticities, ,n, and rj2, are
affected by the same variables as demand.
The member (Argentinian) firm's objective function and first order conditions may
be written similarly, except in that demand in RoW depends explicitly on both Argentina
and non-member prices, with the latter being treated as exogenous.
M4ax' elpx'X) (2)
Max* p;x;(p,,P p;Q,,Y,)+ e2. P2X2-(P29P21Q21Y2)-C*(Xl`)W' -C2(X)' 2
F.O.C.s P;t + . ]- , c;(x;(P1,p;,Q,, Y)) = 0 *P. I Pi (2a)
10 Argentina's price is effectively rolled into the general price level in the rest of the world, captured by the
world's price deflator Q2. The assumption is not unreasonable. Argentina's share of Brazil's imports
exceeds 5% in 22.6% of all HS-6 headings, but in only 3.1% of headings in RoW even using our limited set
of exporters to define world sales.
11
I *T *6i ;
s[ iw 2* ___(2b)
p2[+ C21(x2(p2,p,Q2,YD))=O ; = 17
The first order conditions imply that, for any market and supplier, an increase [n
either the tariff or the supplying country's exogenous wage, or a decrease in the exchange
rate will increase the marginal cost of delivering exports. The supplying firm must
therefore increase its marginal revenue by altering its landed price (p). We have shown in
Appendix I, that the nature of this change depends on how the price elasticity of demand
changes as costs change.
By assuming that the two markets are segmented and have independent cost
functions we are making them strategically separable, so that we can develop two separate
pairs of price equations." In Brazil:
PI =f,(-,Pi,Q1,}1) (1a)
P. = Y.(.l,p,zX) (2a)
and in RoW:
P2 = f2 (-, Q2, Y2) (lb)
e2*
P; = A ( W. ,P2, Q2, Y2) (2b)
e;
I There is strong evidence to support that markets are in fact segmented-see for example Knetter (1 989)
and Marston (1990).
12
These equations are homogeneous of degree one in costs, competitor's price, the aggregate
price and nominal income in local currency. Our assumptions imply that firms play an
interactive pricing game in the Brazilian market, solving (la) and (2a) simultaneously,
while in RoW the solution is recursive with (lb) affecting (2b) but not vice versa.
For estimation purposes we log-linearize equations (1) and (2) and estimate reduced
form equations for prices. Thus,
. .
ln P;= A, +,BIlnWl + 61 nW[1t+a,ln Q, +2i,n Yl (3a) '2
e, el
. .
lnpj =A ; + 61 In w +/,B; In , '+a lnQa + I lnh Y1 (3b)
el e,
lnp2 = A2 +±82 ln-+a2 lnQ2 +22 InY2 (4a)
e2
w~~~~~~~~~~~~~~~~~~~~~~~~~~~~
In p; = A; + 52 In-w+,8 .I2n-, + a* In Q2 + X21 n Y2 (4b)
Equations (4a) and (4b) are written without tariffs in the RoW, i.e., without r2 and T2,
because these variables are considered fixed over our sample period, and thus are absorbed
into the constant term.'3 Feenstra (1989) uses a variant of equation (3a) to show that for
US imports of Japanese trucks and cycles, the long-run pass-through of tariffs and
12 In accordance with the symmetry hypothesis we have given the tariff and wage the same coefficients in
these equations, but in our estimations we separate out the tariffs.
13 In fact these rates did actually change a little over time, but much less than in MERCOSUR. In any case,
since we have no data on 'world' tariffs, these variables must either be taken as constant, or absorbed into the
error term as white noise.
13
exchange rates are statistically identical. Essentially, it focused on the m.f.n. effects, P of
the equation, whereas the coefficient of interest in the 'strategic' pricing relevant to PTAs
is 81*. If marginal costs are fixed then the expected sign of 81* depends only on how its
'perceived' price elasticity of demand gets altered from the preferential tariff inducedl
reduction of its rival's price. If the non-member's demand becomes more elastic, then the
optimal response is to reduce price, hence 8,* > O."4 Detailed analysis and interpretations
of the coefficients and comparative statics is relegated to Appendix I.
While (3) and (4) are estimable directly it is intuitively easier and econometrically
more efficient to combine them into a series of relative price equations. Subtracting (3a)
from (3b) generates an equation for the relative prices of member and non-member country
exports to Brazil. Using the homogeneity assumption, i.e., a1, =1-,6, -E, -Al, ancl
a. = 1-f,l -86 - X, we get:
ln PL = A + (51 - WV1 wn l +W(V8-t5; ) 1n +Y(X-A )l (5)15
pI e, , , , Q,
14 Using the framework of Bulow, Geanakoplos, and Klemperer (1985), we say that the strategic interaction
between these rivals' pricing would be 'strategic complements'. This is what one would expect under price
competition. The less likely outcome is also possible: a reduction in the Argentine price can cause the non-
member's demand curve to become less elastic, at least locally, hence making it optimal to raise price. Thus
'strategic substitutability' is also a possibility, though probably rare.
15 If we were willing to assume symmetry between (3a) and (3b) such that B, =,6; = ,a = a , and =
(5) would simplify to a form expressing relative member/non-member pre-tariff prices for a product as a
function of relative costs and the tariff preference margin: In P' = A + (8 - w/e1 + 5 -,6) The
bar over the price denotes pre-tariff prices.
14
Figure 4, summarizes the effect of a preferential tariff shock on the relative prices.
Panel A describes the 'normal' effect of a preferential reduction of tariffs on a trade
partner. The reduction shifts the member's reaction function rf,* to rf2*, less than
proportionately if there is incomplete pass through. If this were all, and the new
equilibrium were M, the partner price and the price relative (p*/p) would have shifted by
no more than the proportionate change in the tariff factor T*. But, in fact, non-partner
exporters react to the price change, ultimately shifting equilibrium to N. Here both prices
have fallen but the price ratio has fallen by less than at M, and hence certainly less than
proportionately to the tariff shock. In terms of equation (5) the elasticity (I3-o6*) lies
between 0 and 1. It is also possible to have cases such as panel B, where a very responsive
member reaction function causes the elasticity to be greater than 1, and panel C, in which a
very responsive non-member implies a negative elasticity. We have shown that the cost
elasticities can have a wide range, but it is also clear that in all three panels the non-
member price falls. To measure this effect directly we need to isolate 8,*.
Turning to the non-members' equations (3a) and (4a) we can compare relative
export prices to Brazil and RoW. Applying homogeneity again,
Pi / Q, w~~~~ 1i [vi
ln_ -c_ 811 . __ T 2n'~~2iQ =c/ln - 1-0f21n[ +,51In * +A In1 L In (6)
P2 / Q2 [eQlQ |e2QI e1Ql ] l] Q2
Similarly equations (3b) and (4b) for Argentina imply
In__Q__ *+A w~z1 *I I *, WV i wi. Y .
P;IQ2 4 LelQ, A eljQe2Q2J 1 eQ, 2 e2Q2J QQ2 (7)
15
In summary, while equation (5) shows how much the non-member's export price changes
in Brazil relative the member's, export price, equation (6) shows how much the non--
member export price changes relative to non-member exports to RoW, and (7) how much
the member export price changes relative to its export prices to RoW. Our interest is
primarily on how the tariff preferences inherent in MERCOSUR have changed Argentinian
and non-member export prices--i.e. on the coefficients on t, in these equations. Figures 1
and 2 suggest that there were significant effects through time and (5)-(7) help as to identify
whether those are due to tariff changes (MERCOSUR) or to other factors such as exchange
rates or costs.
3. EMPIRICAL IMPLEMENTATION
3.1 MERCOSUR Tariff Policy
MERCOSUR (Mercado Comuxn del Sur) was established under the Treaty oi
Asunci6n, signed by the Presidents of Argentina, Brazil, Paraguay and Uruguay in 26
March 1991 and ratified on 29 November 1991. This treaty extended the borders of the
association between Argentina and Brazil dating from 1985 and culminating in The Treaty
of Integration, Co-operation and Development of November 1988.16
16 Nogues and Quintanilla (1993) note that regional integration efforts between Argentina and Brazil did not
go beyond 'declarative' statements until the Protocols initiated between 1985-1989 on capital goods which
was mainly designed to substitute imports from cheaper sources.
16
Article 5 of the Treaty of Asunci6n defined a path of tariff liberalization to achieve
zero internal tariffs and the elimination of non-tariff barriers by the end of 1994. The
immediate reduction of the internal applied tariff rates was by 47% of the m.fn. rate after
the ratification of the Treaty on 29 November 1991. Subsequent preferential reductions
relative to prevailing m.f.n. rates were to occur semi-annually and automatically according
to the following time table: 54% December 1991, 61% June 1992, 68% December 1992,
75% June 1993, 82% December 1993, 89% June 1994, and finally 100% December 1994.'7
Members were allowed to declare upto 300 exceptions to internal free trade, but by 1995
approximately 95% of intra-regional trade was duty-free--Laird (1997). In fact Brazil had
only 27 exceptions and so effectively had open borders for its MERCOSUR partners.
MERCOSUR member countries had originally planned to align their external
tariffs on the MERCOSUR common external tariff by 1 January 1995. However, this
proved politically impossible and little progress was made in defining the CET until the
Protocol of Ouro Preto was signed in December 1994. Under the Ouro Preto Protocol the
CET was to be introduced beginning 1995. Each member was again allowed an exceptions
list, the tariffs on which were to be aligned by 2001 for Argentina and Brazil, and 2006 for
Paraguay and Uruguay, see Olarreaga and Soloaga (1998). Brazil named approximately
200 tariff lines in the exceptions list, mainly sensitive industries such as computers,
electronics, chemical, agroindustry, textiles, capital goods (machinery), and the automotive
industry. Unilateral liberalization followed by this negotiated changes reduced tariffs
7 Article 3, Annex 1, Trade Liberalization Program, Treaty of Asunci6n, 1991.
17
substantially in MERCOSUR countries, from an average of 50% in 1988 to a CET average
of 12% in 1995. However, it remained the case that trade policy in Brazil was subject to
vigorous debate and to frequent changes to meet short-run political objectives. For
example, tariffs on textiles, toys and motor vehicles in particular were increased to 70% for
non-members in 1995.18
The different phasing of internal and external tariff reductions, the large number of
tariff rates and the use of exceptions mean that over 1989-96--our sample period-tariffs
and preference margins varied widely over time and commodities. This allows us a good
chance of identifying their effects empirically.
3.2 Data
Our trade data, used to obtain unit values from quantities and values, were taken
from the UN's Comtrade database, at the Harmonized System (HS) 6-digit level. Although
it was introduced in 1989 several countries did not start to use HS until somewhat later.
Hence our sample periods vary by country.
HS 6-digit data offer two major advantages over other sources. First, they are very
disaggregated--over 5,000 commodities are distinguished. This helps to minimize
heterogeneity within each heading, which in turn improves the quality of our unit value
" Motor vehicles have been a special issue within Brazil. The Brazilian government applied special local
content rules. Foreign multi-national fiirms which produced vehicles locally were given reduced rates of
35%. Japanese and Korean auto manufacturers in particular claimed that the moves put them at a
considerable disadvantage since, not having local plants, they were not able to compete even with other non-
member suppliers. These types of local content rules prompted several multi-nationals to set up automobile
18
data, and reduces the need for tariff averaging within headings-see next paragraph.
Second, trade and tariff data match very well at the 6-digit level, because at this level the
HS classification is universal across countries. At finer levels of disaggregation codes are
country-specific."9
The tariff data were provided by UNCTAD and the MERCOSUR Secretariat-to
whom we are grateful. Over the years 1989-1994 Brazil and Argentina defined their tariff
data at HS 10-digits, while the Common External Tariff (CET) of 1995 and 1996, and the
exceptions listed in the agreement of Ouro Preto Protocol, are defined at the HS-8 digit
level. In order to concord the tariff and the price data we truncated the tariff codes up to
the 6-digits and took simple averages. This averaging within the HS-6 level is not a
serious problem because there is very little variation in tariffs within the HS-6 digit level.
As an empirical exercise on the price effects of integration, a study of MERCOSUR
is relatively problem-free. There are few problems of changes in quotas confounding price
movements, since on signing of the Treaty of Asunci6n, all non-tariff barriers were to be
removed for all trade including imports from non-members.2" Products having NTB
measures before integration which could potentially affect prices over the series were
plants within the MERCOSUR region. For details see Latin American Monitor-Brazil and Latin American
Regional Report-Brazil, August (1996).
'9 There is a slight discrepancy between the HS-6 digit codes in HS92 and HS96. Commodities have been
deleted when such concordance problems arise between years.
20 See Laird (1997) and Frischtak, Leipziger, Normand (1996). The abolition was not entirely clean in
practice, however. There are some instances where quotas may have been used, particularly in textiles. Due
to heavy losses and high unemployment in the Brazilian textile industry there was great pressure to impose
quotas and high duties, especially against Southeast Asian countries. Quota protection and local content
rules were threatened by Brazil in the automobile industry as a means to attract foreign direct investment, but
19
deleted from our sample altogether.2' Applied tariff rates are entirely ad valorem charged
on the c.i.f. value of imports. There were no major prior associations between these
countries and therefore changes in tariff preferences are defined by the Treaty of Asuncian
and the Ouro Preto Protocol. The first shock comes at the beginning of the transition
period at the very end of 1991, and the effects can be seen in 1992, and 1993. Then
another major shock comes in 1995, when the CET is implemented with exceptions which
tend to increase tariffs on non-members.22
Internal tariff rates were calculated as the m.f.n. rate multiplied by (1 - average
reduction rate for that year). Since the reductions take place semi-annually (see above) we
have to average them for each year to match the annual trade data. The following chart
provides a typical transition for most commodities, although we have incorporated the
exclusions to this rule included in the agreement of Ouro Preto Protocol in December 1994,
which took effect in 1995, as well as the changes that occurred subsequent to this
Protocol.23
after further negotiations with Argentina they were revised and ceased to be binding--see Latin American
Monitor: Southern Cone Report, February 1996.
21 This list, obtained from UNCTAD, includes products under quantity control measures such as quotas, and
voluntary export restraints.
22 Most of the applied m.f.n. tariff rates charged to non-members including exceptions were compiled by
UNCTAD. We are grateful to Aki Kuwahara of UNCTAD and Jerzy Rosanski of the World Bank for their
help in obtaining them. Detailed information can be obtained in United Nations Conference on Trade and
Development (UNCTAD) "A User's Manual for TRAINS", 1996. The internal tariff rates are estimated
using these m.f.n. rates and the Treaty of Asunci6n's time path. Brazil's detailed import and export data
disaggregated by source country were also provided by Aki Kuwahara. Argentina's trade data, which was
used in the intermediate stages of our research, was provided by Tony Estevadeordal and Raphael Comejo of
the Inter-American Development Bank to whom we are also grateful.
23 This list was provided by the MERCOSUR Secretariat.
20
m.f.n rate Internal rate
t89 t89
t9o t9o
t91 t91
t92 t92*(1-0.61 )
t93 t93*(1-0.75)
t94 t94Z(1e .89)
t95 Zero
t96 Zero
As an illustration of the evolution of tariffs, we have tabulated the tariffs charged to USA
(m.f.n.) and Argentina (partner) and the preference margin in Table 1.24 These are HS 6-
digit tariffs truncated up to 2-digits and then averaged (unweighted) across the nine
categories specified in Appendix II. Some notable features are evident even at this
aggregated level. First, although the m.f.n. rates are generally falling after 1991, there are
also some increases in 1995 and 1996 as a result of Ouro Preto--in HS Chapters 16-27
(prepared foodstuffs), 41-63 (which includes textiles), 64-83 (which includes footwear,
headgear, glass etc.,) 86-89 (which includes vehicles, aircraft, vessels, transportation
equipment, etc.) and 93-96 (which includes toys). The increases in 1995 and 1996 were
within Brazil's overall binding commitments at the WTO.
Second, while m.f.n. rates decline from 1991 to approximately 1994 and then
stabilize or rise, the tariffs on partners continue to fall until 1995. Thus member and non-
member tariffs are not perfectly correlated, which greatly facilitates the identification of
4 This table is confirmed by Laird (1997), but unlike Laird, who averages all tariff data available, we
provide the average tariffs only for the commodities for which US export price data are available over the
years 1991-1996, since these are the tariff rates used in the estimation for USA export pricing behavior in the
following section.
21
separate effects econometrically. Third, preference margins did not rise monotonically as
MERCOSUR was implemented.
Finally, member and non-member wage rates or labor costs could not be obtained
at the industry level and certainly not at the commodity level over the time perioil
necessary in this analysis. Thus in order to obtain data and also to recognize a wider range
of inputs than just labor, we used GDP deflators to proxy export country costs (using
aggregate export weights to Brazil to construct non-member costs). These variables could
easily be converted into the currency of the importer.25 For the aggregate price index in
Brazil and RoW we employed GDP deflators.
4. RESULTS
4.1 (A) Relative Import Prices in Brazil
Our main results appear in Tables 2 through 6. As well as pooling all commodities,
these also consider 9 sub-groups of commodities. The disaggregation allows scope for
some variability in the degrees of competition and product substitutability (differentiation)
across sectors. In every panel all variables are expressed in natural logs and as deviations
from commodity-specific means. This is equivalent to allowing commodity-specific fixed
effects. We also corrected for heteroskedasticity by collecting the residuals from the
25 The GDP deflator for the world in dollar terms was taken to be an export weighted average of the GD:P
deflators of supplying countries, with weights coming from the International Monetary Fund, Direction of
Trade Statistics: Yearbook (1996, 1997). The representative countries included in the weighted average are:
22
estimated unweighted equations and reweighting each of the variables by the inverse of the
estimated commodity-specific residual standard deviations.26 This procedure improves the
efficiency of our estimates and permits more accurate inference.
First we examine the prices of Brazil's imports from Argentina relative to a series
of non-member countries, equation (5).27 To try to isolate the effects of most interest, we
have separated out the tariff effects.28 These initial estimates appeared to suffer very
seriously from multicollinearity. This seemed traceable to the coefficients of the real
income terms (Y/Q), which regularly had variance inflation factors above 20 and
frequently much higher. The problem is three-fold. First, Brazil's measured real income
was rather stable over 1989-96 so that there was little identifying power in the series.
Second, with inflation reaching 2308 % in 1994, it was unclear whether deflated nominal
income is really very informative anyway. Third, all the explanatory data except tariffs
refer to macroeconomic variables (the exchange rate, costs, aggregate prices and incomes)
which are invariant over commodities. Thus in effect we are seeking to identify three
effects with eight observations.
Belgium, Bolivia, Canada, Chile, China, Colombia, Denmark, France, England, Germany, Indonesia, Italy,
Korea, Mexico, Malaysia, Netherlands, Peru, Philippines, Singapore, USA, Venezuela.
26 The homoskedasticity assumption was tested by using the log-likelihood ratio test and the null was always
strongly rejected. The procedure adopted is a two step Feasible Generalized Least Squares (FGLS)
estimation, which is unbiased. The coefficient estimates in the first stage regressions were quite similar to
the cross commodity heteroskedasticity corrected set and can be obtained from the authors on request. The
uncorrected estimations tended to yield very low R-squares, however.
27 Brazil is used as the reporter country for the data used in Table 2A and 2B, and therefore the data run from
1989-1996, with the exception of Germany which Brazil only reports from 1991-1996. The countries
represented in Table 2 make up most of the imports to the Brazilian market.
23
We have adopted two approaches to the multicollinearity problem. In estimate (A)
we have assumed that 2A = *,V and dropped the real income term. Strictly this implies that
for each good, the Argentinian and non-member varieties have the same income elasticities
of demand, but it is better thought of as merely as indicating that we have insufficient
information to identify different elasticities. In estimate (B) we have swept out the
macroeconomic effects with time dummies for each year, leaving the tariff effects as the
only explanatory variables. Essentially relative Argentinian and non-member prices
comprise a time-related component, which we isolate and ignore in these equations, and a
commodity-specific component related to the two tariff rates. With some exceptions, the
estimates of the tariff effects--our variables of interest--are similar between the two
approaches.
Tables 2(A) and 2(B) report the results from the overall pooled samples. They
display a number of interesting features. First, tariffs matter for firms' pricing decisions.
Both member and non-member tariffs are strongly statistically significant in explaining the
relative prices of imports within the Brazilian market. Nearly all of the overall results are
highly significant, have the correct signs and have reasonable magnitudes according to our
discussion above.
Second, Brazil's tariff factor on Argentinian imports (T*) affects relative
member/non-member prices less than proportionately in ten out of the twelve cases. With
the exception of Mexico and Japan, the member's tariff coefficients are less than one in
2S The results of equation (5) with the tariffs combined with the rest of costs are shown in the Appendix,
Table Al.
24
Table 2A and not significantly above in Table 2B. The remaining estimates range from
0.282 for Korea to 0.884 for France, and all are statistically significantly different from
one. These latter results reflect some convex combination of (a) Argentinian firms passing
only part of the tariff cut onto consumers (partial pass-through) and non-members holding
their prices constant (8o*=0), and (b) Argentinian firms passing the tariff cut through fully
(P,*=1) and non-member firms partially following iheir prices down (0<68*